Abstract
Political tolerance is a key democratic value believed to undergird successful and healthy democracies. In nascent democracies especially, citizens must tolerate the views and participation of opposing groups in order to ensure methodical transfers of power with successive elections. Yet, despite its importance, little research considers tolerance outside established democracies. In this paper, we compare political tolerance across eight Eastern European countries and six Western countries. We demonstrate that mean levels of tolerance are lower in the newly democratized countries of Eastern Europe and then examine whether they are a function of East Europeans’ limited experience with democracy. We also test whether established individual-level theories of tolerance replicate across this wide range of new and old democracies. We find some support for theories of democratic learning and also show that models of tolerance operate differently across the range of countries in our sample.
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Notes
These studies often include a combination of social, ethnic, and/or political tolerance. For example, Sniderman et al. (2000) and Bahry et al. (1997) explore social tolerance at times in conjunction with political tolerance. Studies of ethnic tolerance are concerned with prejudical attitudes, such as stereotyping, rather than the extension of civil liberties to groups. Studies of ethnic tolerance in Eastern Europe include, for example, Bulgaria and Romania (McIntosh, MacIver, Abele, & Nolle, 1995) and the former Yugoslavia (Hodson, Sekulic, & Massey, 1994; Kunovich & Hodson, 1999). Other studies noted in the overview center specifically on political tolerance.
While general support for democratic values would be expected to increase in a population simply due to the excitement of transition, individuals must ultimately learn the difficult lessons of applying democratic values in practice. Civic education programs are one important place for individuals to learn to move beyond simple “lip service” to true expression of democratic values. International donor agencies were quick to provide democracy assistance of many types, including civic education programs, after the transitions to democracy in Central and Eastern European countries. Each of the Central and Eastern European countries in the analyses (except former East Germany) has been the recipient of democracy assistance from United States Agency for International Development (USAID), receiving aid for between 4 and 14 years (Finkel, Perez-Linan, & Seligson, 2006). The presence and content of the programs is designed to impart the norms of tolerance through the school system, so that with successive groups of individuals socialized under these civic education programs, tolerance may be more diffused across the general publics of former communist countries. And, early program evaluations suggest that students in some former communist countries are embracing civic skills and knowledge imparted through these endeavors (Amadeo, Torney-Purta, Lehmann, Husfeldt, & Nikolova, 2002; Torney-Purta, Lehman, Oswald, & Schulz, 2001).
The relationship between education and tolerance is debated. Some researchers have argued that any relationship between education and tolerance is due to social desirability pressures among the well-educated (e.g., Jackman, 1973). In addition, education may create only superficial support for tolerance (Jackman, 1978; Jackman & Muha, 1984). That is, highly educated individuals may express general tolerance but not support specific policies. Because the political tolerance questions to be used in this analysis are specific, they may avoid the latter critique.
The inculcation of tolerance through democratic activism may also occur because democratic activism is often undertaken by political minorities to change the policies of the government or political majority. Opposing one’s government helps a citizen appreciate the value of procedural rights (Peffley & Rohrschneider, 2003, p. 246). An increased appreciation of rights should translate into increased acceptance for the extension of rights to more unpopular groups.
The list of least-liked groups varied in a few of the countries, reflecting national differences in the relevance of different groups. For example, in the United States respondents were asked about: immigrants without visas, right-wing extremists, Islamic fundamentalists, welfare recipients, criminals, homosexuals, and left-wing extremists. While criminals were included in the original question, analyses presented here exclude individuals who picked criminals as it potentially becomes the default category. This is consistent with previous research (Guerin et al., 2004; Peffley & Rohrschneider, 2003). Results are essentially the same if this category is included. Information about these differences across countries or about the confirmatory factor analyses are available from the authors.
The other approach to measurement is a group-specific approach. Group-specific tolerance research asks questions about particular groups such as communists, socialists, and atheists (Stouffer, 1955; Nunn et al., 1978). Most cross-national research on tolerance has used group-specific measures. For example, Karpov (1999a, b) asks about communists, socialists, and atheists. Katnik (2002) used questions about revolutionaries, while Duch and Gibson’s (1992) work investigated fascists.
Some argue that such democratic actions may be undertaken by a minority to alter, for example, majority influence or existing policies (Peffley & Rohrschneider, 2003; Rohrschneider, 1999). By aggregating these five activities, democratic activism is considered in broad theoretical terms and includes a range of activities that illustrate engagement with the democratic process. We thank the editors for this point.
There are two things to note about our coding. First, by delineating individuals based on receipt of even a single year of education in the democratic or thawed time periods, we are making a conservative coding decision. Second, we tried a variety of alternative codings, including spending “formative years” (through age 25) under a particular regime, and 10-year age cohorts. The results are similar across alternative codings.
All survey respondents included in our analyses reported attending school. Nearly 80% of respondents in Eastern European countries reported at least 10 years of schooling.
While including multiple levels of analysis in cross-national research is useful, we consider it to be a topic beyond the scope of this analysis. Theoretically, a key purpose of our research is investigating individual-level theories of tolerance in countries for which such data are newly available, especially former communist countries in Central and Eastern Europe. Statistically, in this set of countries the across-country variation is very small, with an intraclass correlation coefficient (ICC) of only 3.2%. Empirically, therefore, the influence of any country-level effects that could be assessed would be explaining only a portion of this total variation of around 3%. The multilevel intra-class correlation coefficient in our sample suggests that there are important within-country differences in these two regions that should be taken into account before moving to an examination of other forces (e.g., between-country variation). We argue that it is important to first establish whether the individual-level models of tolerance developed and tested in the West apply in newer democracies in Eastern Europe before seeking country-level explanations for their influence. Including fixed country effects helps account for any unobserved heterogeneity across countries (e.g., in GDP or other national-level confounds). Our results therefore provide one snapshot of how general publics in these countries compare with one another, relative to West Germany for Western European countries and relative to Poland for countries in Central and Eastern Europe.
With such a small sample (eight countries in Eastern Europe and six in the west), we likely do not meet the assumptions of a two-sample t-test, e.g., assumptions of normality and equal variances, that tend to be minimized in large samples. Instead, testing for differences in medians across the regions requires a nonparametric statistic. We use the Fligner–Policello test (Fligner & Policello, 1981; Hollander & Wolfe, 1999) because it does not assume normality or equal variances and, moreover, does not assume similarity in the shape of the distributions. We thank Paul von Hippel for writing a SAS macro to perform the test.
Standard deviations across individuals are also lower within each country of Eastern Europe, when compared to Western Europe and the United States.
Our models shown in Table 2 and Appendix 2 include both age in years and dummy variables for two theoretically derived age cohorts, those experiencing some education under the thawed period and those with no education under the thawed or free period. If we were only to include the age cohorts in the model, we would confound the effects of age and cohort. That is, the hypothesis of increasing conservatism with age must be distinguished from the age group effects. While a model with both age and cohort would traditionally be considered unidentified, when the effect of age is linear and that of cohort nonlinear, the two effects can be distinguished (Kritzer, 1983). However, we must be aware that the introduction of both effects simultaneously may increase multicollinearity. As a robustness check, we ran the model without age and came to similar conclusions. Although the coefficients for the age cohorts change (as would be expected since they are picking up some linear age effects), the coefficients in the West remain nonsignificant and those in the East significant.
Recent reviews of methods to deal with missing data list maximum likelihood as one of two optimal procedures (e.g., Allison, 2002). We used the maximum likelihood missing data procedure in AMOS, a structural equation modeling software package.
Since models of tolerance were mainly developed in studies of the United States, we consider the model of tolerance in that country alone.
Based on prior research, West Germany is the reference category for the model of Western European countries and Poland is the reference category for the model that includes Eastern European countries.
In contrast to Western Europe, with effect sizes that are essentially zero, the coefficients for the United States are increasingly negative, although they cannot be statistically distinguished from zero.
Although the effect size is small, this result is supported by some previous research. In a similar analysis, Duch and Gibson (1992) find differences between age cohorts socialized during nascent democratization in the newer democracies of Greece, Spain, and Portugal.
Although they are often lower than the United States, the coefficients in Western and Eastern Europe do differ significantly from each other in a number of cases. Statistically significant differences between east and west (at conventional levels) occur for the coefficients of the following variables: urban residence, group memberships, female, espoused support for democracy, trust, protectionism, church attendance, and both age cohorts.
In auxiliary analyses we tested additional variables considered in some previous research, such as conformity and government satisfaction. The addition of these variables to the model did not change the basic results as presented here.
This finding is not entirely unexpected, considering theory that suggests that association memberships are not always beneficial to democracy and democratic values, e.g., Paxton (2002).
An auxiliary analysis of the model in Australia produces results similar to those in the United States. Thus, Western Europe may be the exceptional case in its history of warfare and occupation by totalitarian regimes. Future research should consider a wider range of countries, including developing countries.
Even within a region, differences in measurement can lead to differences in results. For example, the model results for Sweden and Finland differ to some degree, which may be a function of differences in the least-liked group options. In Sweden, religious groups were included as an option but not in Finland. This small difference in measurement, even within the same region, could account for the differences seen for the variables church attendance, etc. in Appendix 2.
A limitation of our research, indeed of much research in the area, is the lack of a “standard” specification of tolerance. For example, two previous studies have used the World Values Survey to model tolerance. But the two studies use different sets of variables to predict tolerance, and each of these sets of variables differs from our model. Without a standard model of tolerance, comparisons across contexts will remain difficult. Hopefully, as more research considers which concepts “travel” across countries, we will be able to develop a standard model of tolerance than can be replicated precisely in a variety of contexts.
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Acknowledgments
We thank Bob Kaufman, Lisa Morrison, and Kazimierz Slomczynski for comments on an earlier version of this paper. We gratefully acknowledge the support of the Mershon Center at The Ohio State University. This paper was presented at the 2002 American Sociological Association meetings in Chicago, Illinois.
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Sandra Marquart-Pyatt and Pamela Paxton contributed equally to this paper.
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Marquart-Pyatt, S., Paxton, P. In Principle and in Practice: Learning Political Tolerance in Eastern and Western Europe. Polit Behav 29, 89–113 (2007). https://doi.org/10.1007/s11109-006-9017-2
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DOI: https://doi.org/10.1007/s11109-006-9017-2